The biosphere models of terrestrial productivity are essential for projecting climate change and assessing mitigation and adaptation options. Many of them have been developed in connection to the International Geosphere-Biosphere Program (IGBP) that backs the work of the Intergovernmental Panel on Climate Change (IPCC). In the end of 1990s, IGBP sponsored release of a data set summarizing the model outputs and setting certain norms for estimates of terrestrial productivity. Since a number of new models and new versions of old models were developed during the past decade, these normative data require updating.-1, the productivity of larch forest zone may be comparable to the productivity of taiga zone, and the productivity of rain-green forest zone may be comparable to the productivity of tropical rainforest zone.Here, we provide the series of updates that reflects evolution of biosphere models and demonstrates evolutional stability of the global and regional estimates of terrestrial productivity. Most of them fit well the long-living Miami model. At the same time we call attention to the emerging alternative: the global potential for net primary production of biomass may be as high as 70 PgC yThe departure from Miami model's worldview mentioned above cannot be simply ignored. It requires thorough examination using modern observational tools and techniques for model-data fusion. Stability of normative knowledge is not its ultimate goal – the norms for estimates of terrestrial productivity must be evidence-based. It shows biosphere potential to supply primary food energy source for non-autotrophic species including humans. Human appropriation of terrestrial net primary production stems not only from the demand for food but also for fuel, construction materials, and paper. It is estimated to be from 8 to 15 PgC y2, a gas whose atmospheric concentration affects global climate. NPP characterizes the "gross" terrestrial carbon sink – the amount of CO2 annually sequestered by vegetation. The net land-to-atmosphere flux is much smaller because the "gross" sink is compensated for by various carbon sources. Its magnitude is estimated to be from 0.3 to 1.5 PgC y-1 [-1 [NPP also shows biosphere potential to steer the Earth system by absorbing COAppropriation (or re-direction) of NPP is also one method of climate change mitigation. Protecting non-living organic matter from decomposition and burning , reducin-1 [-1 [The total terrestrial NPP is generally assumed to be about 60 PgC y-1 . Biosphe-1 [-1 . One of -1 [-1 . These wThe data have not been updated since then, although a number of new models and new versions of old models were developed during the last decade. Here we present the series of updates reflecting the evolution of biosphere models.-1. Sub-totals characterizing productivity of major vegetation zones are never off by more than 7 gC m-2 y-1  works against estimates suggesting too large shifts in mean values. These estimates form the pool of alternative data. The series of alternative data . Then, the data was turned into empirical models that relate gradations in NPP to environmental factors of known geographic distribution. Later, a number of process-based models were developed in connection to the IGBP activities. This is definitely a field of science that hardly may be referred to as immature.-1 [-1. The estimates of empirical models [-1, and the estimates of process-based models are expected to vary in the same range [-1 [Nevertheless, the range of estimates remains roughly constant over this period. Early estimates of terrestrial NPP range from 10 to 100 PgC y-1 . Startinl models  vary froange [-1 . It seem-1 remains to be the best candidate for further use in policy relevant assessments.Therefore, we are focusing here on the stability of normative estimates – that is, estimates acceptable for use in policy relevant assessments. The diversity of research results does not matter until a viable alternative to the commonly accepted norms emerges. This study confirms that 60 PgC yThe major output of this study, gridded normative data on terrestrial productivity, may find use in benchmarking NPP models employed in coupled carbon-cycle-climate models. Could the wide range of projections for the future magnitude of the terrestrial uptake be attributed to the diversity of NPP models employed? Which projections correspond to well-established beliefs, and which do not?Recent IPCC guidelines focus on objective reporting of uncertainty stemming from climate model pluralism . HoweverObjective reporting of well-established beliefs suggests drawing distinctions between normative knowledge (or text-book knowledge) and alternative knowledge (or frontier knowledge). The former is the solid knowledge that has stood the test of time and is well confirmed by a number of independent research studies. Frontier knowledge is something new, and something really new cannot be turned into solid knowledge immediately. Since each model may be considered as normative for some regions and as alternative for other regions, we are not drawing distinctions between models. Instead, we are sorting model outputs and doing what is called "knowledge engineering" .2 fluxes offers some benefits over traditional methods of measuring productivity. The observation network for measurements of gas, water and energy exchange between terrestrial ecosystems and the atmosphere, so-called FLUXNET [The stability of well-established beliefs stems from the stability of research methods, and therefore it can be temporal. Rapid development of meteorological methods for measuring CO FLUXNET , produce FLUXNET ,31, and,Modern observational tools will eventually improve the consistency of biosphere models through creating multiple constraints for positioning 'true' values for model parameters -35 and tThe evolution of scientific theories is often considered a Darwinian process of natural selection that determines which theory survives and drifts them toward consensus . The schδ = 2 c·s·n-1/2, where n is the number of estimates, s is standard deviation, c is 95th percentile of Student's t distribution with n-1 degree of freedom. All this is filtering out erroneous estimates as well as correct estimates when they are dramatically contradicting normative knowledge formed by the initial ensemble.This algorithm works against new estimates that do not fall within the range implied by the initial ensemble: Miami NPP model, Montreal NPP model, TGER-NPP model and the outputs of the Potsdam NPP model intercomparison. Moreover, it works against new estimates that may increase uncertainty in the mean value, which is measured as the width of the confidence interval: Since no well-agreed-upon method exists at the moment for distinguishing between erroneous and correct estimates of NPP, every estimate is included either into the normative ensemble or into the alternative ensemble. The former represents the current state of knowledge, whereas the latter represents emerging alternatives to current knowledge.Noteworthy also are discrepancies in estimates that may result from different spatial resolution of models and/or input data. They were reduced by filtering out models of low spatial resolution.This data file  of the geographic grid, u with the normative ensemble of estimates for this cell, w.max or u < wmin, normative NPP, v, remains unchanged; otherwise go to step 3.2. If u > wμ, of thus formed list of estimates and the width of its confidence interval, δ.3. Append u to w, calculate mean value, δ is greater than the width of confidence interval for the mean value of w, v remains unchanged, otherwise v = μ.4. If NB. Exclude the numbers denoting missing values (-9999) from calculations of mean values and their confidence intervals.Each of the data files  of the geographic grid, u with the normative ensemble of estimates (version 1.5.0) for this cell, w.min < u < wmax, alternative NPP, v, remains unchanged; otherwise go to step 3.2. If wδ.3. Append u to w, calculate mean value of thus formed list of estimates and the width of its confidence interval, δ is less than the width of confidence interval for the mean value of w, v remains unchanged, otherwise go to the step 5.4. If μ, of thus formed list; set v = μ.5. Append u to the list of alternative estimates, s; calculate the mean value, NB. Exclude the numbers denoting missing values (-9999) from calculations of mean values and their confidence intervals.Each of the data file  developing an empirical model interpolating and extrapolating measured NPP values, (type 2) developing an empirical model interpolating and extrapolating parameters of a process-based model of productivity using measured NPP values, (type 3) developing an empirical model interpolating and extrapolating parameters of a process-based model of productivity using measured values of this parameters.The typical representative of the first type, which is referred to as empirical models, is the Miami NPP model. The observed gradations in the observed data are attributed to factors of known global distribution using mathematical functions that have no biological meaning. Then, these functions are used to produce a global pattern of productivity from given global patterns of mean annual temperature and precipitations.TsuBiMo represents the second type, which is referred to as semi-empirical process-based models. The measured values of NPP are considered as indirect measurements of light-saturated rates of photosynthesis . The valBiome-BGC represents the third type, which is referred to as process-based models, and GLO-PEM represent a class of so-called production efficiency models (PEMs). A number of such models were developed between 1992 and 1996  together with growing databases of NPP measurements offer new opportunities for empirical modelling. The NCEAS model  may be aThe history of modelling efforts is presented in Table Gridded normative NPP, version 1.5.0. The file can be viewed with a spreadsheet. It stores tabular data: 3 columns separated by tabs and 62483 rows. Each row corresponds to a cell of the geographic grid of half-degree resolution. First two columns contain longitude and latitude for the north-west corner a cell in decimal degrees. West longitudes and south latitudes are negative. The third column contains NPP estimate in gC m-2 y-1 obtained by averaging outputs of Miami NPP model, Montreal NPP model, TGER-NPP model and the outputs of the Potsdam NPP model intercomparison (PotsdamNPP). Missing values are denoted by -9999.Click here for fileGridded normative NPP, version 1.6.0. The file can be viewed with a spreadsheet. It stores tabular data: 3 columns separated by tabs and 62483 rows. Each row corresponds to a cell of the geographic grid of half-degree resolution. First two columns contain longitude and latitude for the north-west corner a cell in decimal degrees. West longitudes and south latitudes are negative. The third column contains NPP estimate in gC m-2 y-1 formed from the normative ensemble of estimates underlying Normative NPP 1.5.0. and outputs of TsuBiMo 1.0 model. Missing values are denoted by -9999.Click here for fileGridded normative NPP, version 1.7.0. The file can be viewed with a spreadsheet. It stores tabular data: 3 columns separated by tabs and 62483 rows. Each row corresponds to a cell of the geographic grid of half-degree resolution. First two columns contain longitude and latitude for the north-west corner a cell in decimal degrees. West longitudes and south latitudes are negative. The third column contains NPP estimate in gC m-2 y-1 formed from the normative ensemble of estimates underlying Normative NPP 1.6.0. and outputs of the updated version of GLO-PEM. Missing values are denoted by -9999.Click here for fileGridded normative NPP, version 1.8.0. The file can be viewed with a spreadsheet. It stores tabular data: 3 columns separated by tabs and 62483 rows. Each row corresponds to a cell of the geographic grid of half-degree resolution. First two columns contain longitude and latitude for the north-west corner a cell in decimal degrees. West longitudes and south latitudes are negative. The third column contains NPP estimate in gC m-2 y-1 formed from the normative ensemble of estimates underlying Normative NPP 1.7.0. and outputs of Biome-BGC 4.1.1. Missing values are denoted by -9999.Click here for fileGridded normative NPP, version 1.9.0. The file can be viewed with a spreadsheet. It stores tabular data: 3 columns separated by tabs and 62483 rows. Each row corresponds to a cell of the geographic grid of half-degree resolution. First two columns contain longitude and latitude for the north-west corner a cell in decimal degrees. West longitudes and south latitudes are negative. The third column contains NPP estimate in gC m-2 y-1 formed from the normative ensemble of estimates underlying Normative NPP 1.8.0. and outputs of BEAMS. Missing values are denoted by -9999.Click here for fileGridded normative NPP, version 1.10.0. The file can be viewed with a spreadsheet. It stores tabular data: 3 columns separated by tabs and 62483 rows. Each row corresponds to a cell of the geographic grid of half-degree resolution. First two columns contain longitude and latitude for the north-west corner a cell in decimal degrees. West longitudes and south latitudes are negative. The third column contains NPP estimate in gC m-2 y-1 formed from the normative ensemble of estimates underlying Normative NPP 1.9.0. and outputs of Madison NPP model. Missing values are denoted by -9999.Click here for fileGridded normative NPP, version 1.11.0. The file can be viewed with a spreadsheet. It stores tabular data: 3 columns separated by tabs and 62483 rows. Each row corresponds to a cell of the geographic grid of half-degree resolution. First two columns contain longitude and latitude for the north-west corner a cell in decimal degrees. West longitudes and south latitudes are negative. The third column contains NPP estimate in gC m-2 y-1 formed from the normative ensemble of estimates underlying Normative NPP 1.10.0. and outputs of Improved MODIS Collection 4.8 NPP. Missing values are denoted by -9999.Click here for fileGridded normative NPP, version 1.12.0. The file can be viewed with a spreadsheet. It stores tabular data: 3 columns separated by tabs and 62483 rows. Each row corresponds to a cell of the geographic grid of half-degree resolution. First two columns contain longitude and latitude for the north-west corner a cell in decimal degrees. West longitudes and south latitudes are negative. The third column contains NPP estimate in gC m-2 y-1 formed from the normative ensemble of estimates underlying Normative NPP 1.11.0. and outputs of Sim-CYCLE(rev). Missing values are denoted by -9999.Click here for fileGridded normative NPP, version 1.13.0. The file can be viewed with a spreadsheet. It stores tabular data: 3 columns separated by tabs and 62483 rows. Each row corresponds to a cell of the geographic grid of half-degree resolution. First two columns contain longitude and latitude for the north-west corner a cell in decimal degrees. West longitudes and south latitudes are negative. The third column contains NPP estimate in gC m-2 y-1 formed from the normative ensemble of estimates underlying Normative NPP 1.12.0. and outputs of VEGAS. Missing values are denoted by -9999.Click here for fileGridded alternative NPP, version 1.6.0. The file can be viewed with a spreadsheet. It stores tabular data: 3 columns separated by tabs and 62483 rows. Each row corresponds to a cell of the geographic grid of half-degree resolution. First two columns contain longitude and latitude for the north-west corner a cell in decimal degrees. West longitudes and south latitudes are negative. The third column contains NPP estimate in gC m-2 y-1 formed from the alternative ensemble of estimates underlying Alternative NPP version 1.5.0. and outputs of TsuBiMo 1.0. Missing values  are denoted by -9999.Click here for fileGridded alternative NPP, version 1.7.0. The file can be viewed with a spreadsheet. It stores tabular data: 3 columns separated by tabs and 62483 rows. Each row corresponds to a cell of the geographic grid of half-degree resolution. First two columns contain longitude and latitude for the north-west corner a cell in decimal degrees. West longitudes and south latitudes are negative. The third column contains NPP estimate in gC m-2 y-1 formed from the alternative ensemble of estimates underlying Alternative NPP version 1.6.0. and outputs of the updated version of GLO-PEM. Missing values  are denoted by -9999.Click here for fileGridded alternative NPP, version 1.8.0. The file can be viewed with a spreadsheet. It stores tabular data: 3 columns separated by tabs and 62483 rows. Each row corresponds to a cell of the geographic grid of half-degree resolution. First two columns contain longitude and latitude for the north-west corner a cell in decimal degrees. West longitudes and south latitudes are negative. The third column contains NPP estimate in gC m-2 y-1 formed from the alternative ensemble of estimates underlying Alternative NPP version 1.7.0. and outputs of Biome-BGC 4.1.1. Missing values  are denoted by -9999.Click here for fileGridded alternative NPP, version 1.9.0. The file can be viewed with a spreadsheet. It stores tabular data: 3 columns separated by tabs and 62483 rows. Each row corresponds to a cell of the geographic grid of half-degree resolution. First two columns contain longitude and latitude for the north-west corner a cell in decimal degrees. West longitudes and south latitudes are negative. The third column contains NPP estimate in gC m-2 y-1 formed from the alternative ensemble of estimates underlying Alternative NPP 1.8.0. and outputs of BEAMS. Missing values are denoted by -9999.Click here for fileGridded alternative NPP, version 1.10.0. The file can be viewed with a spreadsheet. It stores tabular data: 3 columns separated by tabs and 62483 rows. Each row corresponds to a cell of the geographic grid of half-degree resolution. First two columns contain longitude and latitude for the north-west corner a cell in decimal degrees. West longitudes and south latitudes are negative. The third column contains NPP estimate in gC m-2 y-1 formed from the alternative ensemble of estimates underlying Alternative NPP 1.9.0. and outputs of Madison NPP model. Missing values are denoted by -9999.Click here for fileGridded alternative NPP, version 1.11.0. The file can be viewed with a spreadsheet. It stores tabular data: 3 columns separated by tabs and 62483 rows. Each row corresponds to a cell of the geographic grid of half-degree resolution. First two columns contain longitude and latitude for the north-west corner a cell in decimal degrees. West longitudes and south latitudes are negative. The third column contains NPP estimate in gC m-2 y-1 formed from the alternative ensemble of estimates underlying Alternative NPP 1.10.0. and outputs of Improved MODIS Collection 4.8 NPP. Missing values are denoted by -9999.Click here for fileGridded alternative NPP, version 1.12.0. The file can be viewed with a spreadsheet. It stores tabular data: 3 columns separated by tabs and 62483 rows. Each row corresponds to a cell of the geographic grid of half-degree resolution. First two columns contain longitude and latitude for the north-west corner a cell in decimal degrees. West longitudes and south latitudes are negative. The third column contains NPP estimate in gC m-2 y-1 formed from the alternative ensemble of estimates underlying Alternative NPP 1.11.0. and outputs of Sim-CYCLE2. Missing values are denoted by -9999.Click here for fileGridded alternative NPP, version 1.13.0. The file can be viewed with a spreadsheet. It stores tabular data: 3 columns separated by tabs and 62483 rows. Each row corresponds to a cell of the geographic grid of half-degree resolution. First two columns contain longitude and latitude for the north-west corner a cell in decimal degrees. West longitudes and south latitudes are negative. The third column contains NPP estimate in gC m-2 y-1 formed from the alternative ensemble of estimates underlying Alternative NPP 1.12.0. and outputs of VEGAS. Missing values are denoted by -9999.Click here for file


Increased mammographic breast density is a moderate risk factor for breast cancer. Different scales have been proposed for classifying mammographic density. This study sought to assess intra-rater agreement for the most widely used scales  and compare them in terms of classifying mammograms as high- or low-density.The study covered 3572 mammograms drawn from women included in the DDM-Spain study, carried-out in seven Spanish Autonomous Regions. Each mammogram was read by an expert radiologist and classified using the Wolfe, Tabár, BI-RADS and Boyd scales. In addition, 375 mammograms randomly selected were read a second time to estimate intra-rater agreement for each scale using the kappa statistic. Owing to the ordinal nature of the scales, weighted kappa was computed. The entire set of mammograms (3572) was used to calculate agreement among the different scales in classifying high/low-density patterns, with the kappa statistic being computed on a pair-wise basis. High density was defined as follows: percentage of dense tissue greater than 50% for the Boyd, "heterogeneously dense and extremely dense" categories for the BI-RADS, categories P2 and DY for the Wolfe, and categories IV and V for the Tabár scales.There was good agreement between the first and second reading, with weighted kappa values of 0.84 for Wolfe, 0.71 for Tabár, 0.90 for BI-RADS, and 0.92 for Boyd scale. Furthermore, there was substantial agreement among the different scales in classifying high- versus low-density patterns. Agreement was almost perfect between the quantitative scales, Boyd and BI-RADS, and good for those based on the observed pattern, i.e., Tabár and Wolfe (kappa 0.81). Agreement was lower when comparing a pattern-based (Wolfe or Tabár) versus a quantitative-based (BI-RADS or Boyd) scale. Moreover, the Wolfe and Tabár scales classified more mammograms in the high-risk group, 46.61 and 37.32% respectively, while this percentage was lower for the quantitative scales (21.89% for BI-RADS and 21.86% for Boyd).Visual scales of mammographic density show a high reproducibility when appropriate training is provided. Their ability to distinguish between high and low risk render them useful for routine use by breast cancer screening programs. Quantitative-based scales are more specific than pattern-based scales in classifying populations in the high-risk group. Increased mammographic breast density is a moderate independent risk factor for breast cancer, ranking only behind age and family history of breast cancer,2. DiffeFirst edition of BI-RADS classification described four categories of density patterns, but later 2003 edition define them based on quantitative criteria as a quartiles of density percentages ,9. At thThis present study sought to: assess the intra-rater reproducibility of the different visual scales of classification; and compare agreement among the Wolfe, Tabár, BI-RADS and Boyd scales in terms of classification into high- versus low-density groups. Lastly, we explored the variability of the Wolfe and Tabár qualitative pattern-based classifications with respect to the six categories of the Boyd semiquantitative scale.Coruña), Palma de Mallorca, Pamplona, Valencia and Zaragoza from September 2006 through June 2007 were invited to participate in the study. Exclusion criteria of DDM-Spain included the following: women not born in Spain; evidence of previous breast or ovarian cancer; inability to answer the questionnaire; physical impairment to perform the mammogram; and previous breast surgery or implants. The study was reviewed and approved by the Bioethics Committee of the Instituto de Salud Carlos III (Madrid) and all subjects provided written consent. The intended sample size was 500 women per center, implying a total of 3500 women. The final sample consisted of 3572 women (range 496 to 534 per center). The average participation rate was 74.5%, ranging from 64.7% in Corunna to 84.0% in Zaragoza. Mammogram quality at each of these centers had been explored by a pilot study using 25 mammograms per center. As the cranio-caudal (CC) projection posed fewer technical problems, we decided to evaluate mammographic density using the CC projection from the left breast.We used data drawn from the "Determinants of Density in Mammography in Spain" (DDM-Spain) study. This was a cross-sectional study which aimed to identify genetic, reproductive and lifestyle characteristics associated with mammographic patterns/densities that might enhance the risk of developing breast cancer. Briefly, women aged 45 and over who attended the regional Breast Cancer Screening Programs at the recruiting centers established in Barcelona, Burgos, Corunna  and the remaining 1532 were in digital format .Mammographic density was evaluated by a single experienced radiologist . All mammograms were classified using the Wolfe, Tabár, BI-RADS and Boyd scales: analog mammograms were read in negatoscope (view box), and digital mammograms on a computer screen. Every mammogram  was read randomly in four different stages to prevent recall bias until every scale was completed to study comparability among scales.To obtain a subsample of twice-read mammograms to explore intra-observer agreement each of the participant centers chose a set of 50-60 consecutive images using a random number between the first and last mammograms as initial. If the random number proposed was too near the last, the reading continued from the first. The second reading of a total of 375 mammograms was performed with the same procedure than first beginning within 1 to 66 days after the end of the previous reading to prevent recall bias.Intra-observer agreement was evaluated using the kappa coefficient, since it requires no assumption about correct categorization and includes correction for the degree of agreement that would be expected by chance alone. Due to Wolfe scale is compound of four categories defined as N1, P1, P2 and DY while Tabár proposed five different patterns from I to V  and 0.67 , and weighted kappa values were 00.87  and 0.78 , respectivelyComparisons between first and second measures with Tabár classification are shown in Table Kappa values for analog and digital measures were 0.77  and 0.62 , and weighted kappa values were 0.75  and 0.64 , respectively.Using the BI-RADS scale, 315 of 375 images were classified in the same category (84% agreement). No observations differed in more than one category, as can be seen in Table Kappa values for analog and digital measures were 0.76  and 0.76 , and weighted kappa values were 0.90  and 0.90 , respectively.Table Kappa values for analog and digital measures were 0.70  and 0.63 , and weighted kappa values were 0.92  and 0.91 , respectivelyHigh/low-risk categorizations enabled the four scales to be compared. The bivariate study of interscale agreement is summarized in Table Wolfe classified 53.39% of the mammograms as low-risk and 46.61% as high-risk. Total agreement between Wolfe's and the Boyd, BI-RADS and Tabár scales was 75.25%, 75.22% and 90.37% with kappa values of 0.49, 0.49 and 0.80, respectively.Using Tabár's classification, 37.32% mammograms were classified as high-risk.Whereas Tabár displayed almost perfect agreement with Wolfe ), agreement with BI-RADS and Boyd was lower , with kappa values of 0.64 for both (p < 0.0001)BI-RADS classified 21.89% mammograms as high-risk. BI-RADS agreement with the other classifications was as follows: Wolfe, 75.22%, and a kappa value of 0.48 (p < 0.0001); Tabár, 84.46%, and a kappa value of 0.64; BI-RADS and Boyd, almost perfect .Boyd classified 78.14% of mammograms as low-risk and 21.86% as high-risk. It showed almost perfect agreement with BI-RADS , good agreement with Tabár , and moderate agreement with Wolfe .Figure None of the mammograms of low-risk categories of the Wolfe scale, N1 and P1, were included in the high-risk categories of the Boyd scale (E and F) and only one mammogram classified as low risk, category III, using Tabár scale was included in the high risk category E of Boyd scale. Categories N1 (Wolfe) and II (Tabár) are classified mainly into categories A or B of Boyd scale (≤ 10% of density) and categories P1 and III into B, C or D (> 0-50%). On the other hand, 24.75% of the mammograms included in high-risk categories P2 and DY using Wolfe method, were included in low-risk categories, almost all in D category, using Boyd scale. For Tabár scale the percentage was lower, 15.48% (15.34% in D).Category-D mammograms, namely, those with 25% to 50% density, displayed a more heterogeneous distribution under Wolfe's and Tabár's systems.Different methods in use, subjective classifications, and the lack of a gold standard mean that misclassification is a potential problem when assessing mammographic density and potential breast cancer risk. Previous studies have measured inter- and intra-observer variability using experienced radiologists. Most of these studies report good or very good agreement-23. ReliIn our study, good intra-observer reproducibility was observed, particularly when weighted kappa was used. Only the Wolfe and Tabár scales showed disagreement in more than one category in 7 (1.86%) and 16 (4.27%) cases, respectively: six of these cases were classified on both scales with disagreement in two categories, perhaps due to specific characteristics of these mammograms which hindered their evaluation. Only one classification -Boyd's- registered a kappa value of under 0.70, namely, 0.68. Nevertheless, all the disagreements observed (25.06%) corresponded to differences in only one category. It should be noted that Boyd's classification is divided into six categories, with three categories -A, B and C- classifying densities under 25 percent with narrower range intervals than the rest. Half of all mammograms with different results in both readings belonged to these three categories. Taking into account the number of categories and the semi-quantitative nature of the Boyd scale, weighted kappa is a more appropriate estimator of concordance. Using this statistic, concordance for Boyd's scale was 0.92. The other classifications registered agreement percentages of 82% to 84%, and weighted kappa values of over 0.75. Previous studies using the BI-RADS scale reported moderate agreement, with kappa statistics of 0.43 to 0.59 for intra-observer studies. We obtaWhen comparing the different scales, kappa values for distinguishing high-density mammographic patterns ranged from 0.79 to 0.86, revealing almost perfect agreement. This good correlation among the different scales explains the consistency of results on the relationship between mammographic density and breast cancer obtained from different studies using different scales,2. It isSeparate comparison between digital and analog images failed to reveal relevant differences, yielding weighted kappa values for analog versus digital of: 0.87 versus 0.78 using Wolfe's scale; 0.75 versus 0.64 using Tabár's scale; 0.90 versus 0.90 using the BI-RADS scale; and 0.92 versus 0.91 using Boyd's scale. Even though these differences did not attain statistical significance, the kappa values were always slightly higher when our reader examined analog images. This may reflect his longer experience with the old technology, since digital mammographic technology has only recently been introduced in Spanish screening programs.Limitations are intra-observer design of the study and the lack of comparison with computer-assisted methods, which would result in more objective measurements of breast density, slightly higher agreement values, and the possibility of obtaining a measure of density percentages as a continuous variable. Furthermore, such methods are also dependent on observer experience, since the program has to be given some pointers to enable it to delimit the area in which it must calculate the percentage of the breast occupied by dense tissue,2. This Visual classification of mammographic density patterns, besides being quick and easy to perform, is a relatively inexpensive method to implement in breast cancer screening programs. Our study confirms that, using an experienced reader, the four scales display very high reproducibility and are extremely consistent in identifying women with high-density patterns. Quantitative-based scales are more specific in classifying populations in the high-risk group.BI-RADS: Breast Imaging Reporting and Data System; DDM-Spain study: Determinants of Density in Mammography in SpainThe authors declare that they have no competing interests.MP was responsible for the study concept. NA, IGR, CSC, CS, PM, CV & DS were the main responsible of DDM-Study in the corresponding screening programs and provided valuable input to the study desing. FRP & JM conducted a pilot study to verify the quality of mammograms in the different programs. FRP evaluated mammographic density using the four scales. MG & MP performed the statistical analysis and prepared the first draft of the manuscript. The rest of the authors were involved in the field work of the study, recruiting participants, collecting the corresponding mammograms and the information needed to carry out the study. All authors critically revised the manuscript and suggested helpful comments that were subsequently introduced in the text. All authors read & approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2407/10/485/prepub


SMPD1) gene. In Niemann-Pick patients, SMPD1 gene defects are reported to be associated with a severe reduction in plasma high-density lipoprotein (HDL) cholesterol.Niemann-Pick disease type A and B is caused by a deficiency of acid sphingomyelinase due to mutations in the sphingomyelin phosphodiesterase-1  and a hexanucleotide repeat sequence within the signal peptide region, were investigated in 118 unrelated subjects of French Canadian descent with low plasma levels of HDL-cholesterol (< 5th percentile for age and gender-matched subjects). Control subjects (n = 230) had an HDL-cholesterol level > the 25th percentile.Two common coding polymorphisms in the p = 0.317). The frequency of 6 and 7 hexanucleotide repeats was 46.2% and 46.6% respectively in controls, compared to 45.6% and 49.1% in subjects with low HDL-cholesterol (p = 0.619). Ten different haplotypes were observed in cases and controls. Overall haplotype frequencies in cases and controls were not significantly different.For G1522A the frequency of the G and A alleles were 75.2% and 24.8% respectively in controls, compared to 78.6% and 21.4% in subjects with low HDL-cholesterol (SMPD1 gene locus are not associated with low HDL-cholesterol levels in the French Canadian population.These results suggest that the two common coding variants at the  A low plasma level of high-density lipoprotein (HDL) cholesterol is defined as a cardiovascular risk factor and is part of the assessment of global cardiovascular risk stratification . TherapeSMPD1 gene. SMPD1 gene defects are reported to be associated with a severe reduction in plasma HDL-cholesterol  ATP  by standard procedures. Polymerase chain reaction products were resolved on a 6% polyacrylamide denaturating gel.DNA was isolated from the buffy coat obtained after centrifugation of whole blood. Two previously reported common polymorphisms of the  A and B ,15 were G508R) rs50239 wasThe data was analyzed by examining allele frequencies in subjects with low HDL-cholesterol versus controls. The DeFinetti program  was emplHaplotype frequencies (containing both polymorphisms) were estimated for cases and controls, using PHASE version 2.02 .Power calculations using the Genetic Power Calculator  demonstr2 analysis  was performed with respect to allele frequencies in each of the HDL-cholesterol groups. Age, body mass index (BMI), and all lipid parameters in both groups were treated as continuous variables. Comparisons were made through generalized linear model procedures (Proc GLM) followed by Duncan's post hoc test. All p-values < 0.05 were considered significant.Statistical analyses were performed with the SAS package version 8  and SigmaStat version 2.0 . A χth percentile for cases (n = 118), and an HDL-cholesterol > 25th percentile for controls (n = 230). Mean ages were 50 ± 10 and 50 ± 9 years for control and case groups, respectively. Additional demographic and biochemical characteristics are shown in Table p < 0.001). Plasma triglyceride levels were higher in the low HDL-cholesterol group than in the controls .We analyzed a total of 348 subjects from a pool of control subjects and patients with premature CAD. The selection criterion was an HDL-cholesterol < 5ases n = 18, and aSMPD1 gene locus: G1522A (G508R) and a hexanucleotide repeat sequence CTGG (TC)(GT). From our 348 subjects analyzed, 230 controls and 117 cases were successfully genotyped. The prevalence of the G508R variant in cases and controls is shown in Table p = 0.317). The G allele has been reported at a frequency of 0.85 in an European population [p = 0.06, data not shown). Genotype frequencies for the GG, GA and AA classes did not differ between subjects with low HDL-cholesterol and controls (Table p = 0.0003), total cholesterol (p = 0.03) and LDL-cholesterol (p = 0.02) in controls. In addition, we found significant associations between the genotypic classes with total cholesterol (p = 0.009) in the low HDL-cholesterol subjects. We did not find any significant difference in the prevalence of the GG or AG + AA genotypes with the presence of diabetes. The analysis was also carried out separately between each genotypic class , and the results were similar as those presented in Table We examined two polymorphisms at the pulation . We alsop-value of 0.6 after a 1000-simulation analysis. We examined the most frequent genotypes with respect to age, gender, BMI, diabetes mellitus, CAD, family history of CAD, triglycerides, total cholesterol and LDL-cholesterol levels and we found significant differences between the subgroups of subjects with 6/6, 6/7, 7/7 for hypertension (p = 0.04) and triglycerides (p = 0.005) in low HDL-cholesterol subjects only (Table The second polymorphism consisted of a unique hexanucleotide sequence CTGG(TC)(GT) located within the signal peptide region of the acid sphingomyelinase (corresponding to the hydrophobic sequence LVLALALALALA). The genotype distribution of the hexamer polymorphism was examined and the most frequent allele was the 6 and 7 repeats (respectively 46% and 47% of the control group) (Table  p = 0.00 in low Hp = 0.5) , the clinical relevance of these findings is uncertain and are probably the result of multiple statistical tests.Some significant associations were found between th percentile and > 25th percentile for cases and controls, respectively. These data should still be confirmed in a large-scale study.This study is limited by the relatively small number of subjects (n = 348). However, we did have greater than 80% power to detect a genetic variant that accounts for as little as 2% of the variance of HDL-cholesterol. In addition, we used arbitrary cut-points of an HDL-cholesterol < 5SMPD1 locus can cause Niemann-Pick disease types A and B and the concomitant low HDL-cholesterol, the two common coding non-synonymous variants that we examined at this locus do not appear to influence HDL-cholesterol levels to any great extent. Forty-five mutations in SMPD1 gene causing different forms of Niemann-Pick disease type A and B have been described [SMPD1 gene, affecting the activity of acid sphingomyelinase, might contribute to the modulation of HDL-cholesterol levels in the general population. This study did not examine rare mutations and thus carrier status for Niemann-Pick disease type B was not ruled out in either group. However, Niemann-Pick disease type B should not be common enough to influence our findings.Our data suggest that while rare mutations at the escribed . Since tescribed . It remaBMI, body mass index; CAD, coronary artery disease; HDL, high-density lipoprotein; LCAT, lecithin:cholesterol acyltransferase; LDL, low density lipoprotein; SMPD1, sphingomyelin phosphodiesterase-1.The author(s) declare that they have no competing interests.ZD and JCE carried out the analysis of data, participated in the design of the study and the writing of the paper; ILR and ZD carried out the genotyping of the samples; ILR participated in the data analysis and the writing of the paper; MM participated in the patient collection and characterization, the design of the study, the genotyping of the samples, the data analysis and the writing of the paper; JG carried out the clinical examination and collection of the patients, the design of the study and participated in the writing of the paper. All the authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:


The correlation coefficients for patients with benign prostatic disease and prostate cancer were rs=0.632, P<0.001 and rs=0.663, P<0.001, respectively. These results indicate that the relationship between interleukin-6 and C-reactive protein is similar in patients with benign and malignant prostate disease.The relationship between interleukin-6 and C-reactive protein was evaluated in patients with benign ( The systemic inflammatory response is an obligatory response of the body to infection, surgery or trauma. There is also increasing evidence that the presence of a systemic inflammatory response, as evidenced by increased circulating concentrations of C-reactive protein, is associated with early recurrence and poor survival in a variety of common hormone-independent tumours  and in prostate cancer.−1 and/or an abnormal digital rectal examination.Consecutive patients undergoing diagnostic transrectal ultrasound-guided (TRUS) biopsy of prostate were included in the study. The indication for TRUS biopsy was either a total PSA concentration greater than 4 ng mlPrior to biopsy, a blood sample was obtained and the serum stored at −20°C prior to analysis. No patient had a digital rectal examination within a 2-week period prior to sampling. No patient was receiving treatment at the time of blood sampling.None of the patients had participated in a formal screening programme. All patients underwent a minimum of systematic sextant biopsy. A Gleason score was allocated for each tumour. The Gleason score reflects tumour heterogeneity by combining primary and secondary patterns into a composite score, which has been shown to be an important predictor of disease recurrence and survival . Inter-assay variability was <6% for total PSA.−1 and the intra-assay variability was less than 6% over the sample concentration range.Interleukin-6 concentrations were measured using a sensitive solid-phase enzyme-linked immunosorbent assay . The lower level of detection was 0.2 pg l−1 and logarithmic thereafter. Inter-assay variation was less than 10% over the sample concentrations range.C-reactive protein concentration was measured using a sensitive double-antibody sandwich ELISA with rabbit anti-human C-reactive protein and peroxidase-conjugated rabbit anti-human C-reactive protein. The assay was linear from 0.1 to 5 mg lU test. Correlations between two variables were calculated using Spearman's rank correlation test. As the distribution of interleukin-6 and C-reactive protein concentrations was skewed, they were logarithmically transformed to illustrate their relationship in patients with benign prostatic disease and prostate cancer. Analysis was carried out using the statistics package SPSS .Data are presented as the median (range), and, where appropriate, differences between control and cancer groups were examined using the Mann–Whitney P<0.001). There were no significant differences in circulating concentrations of interleukin-6 and C-reactive protein between patients with benign disease and prostate cancer.A total of 145 patients (59 with benign prostatic disease and 86 with prostate cancer) were included in the study. The characteristics of the patients with benign prostatic disease and prostate cancer are shown in rs=0.632, P<0.001 and rs=0.663, P<0.001, respectively.The relationship between circulating concentrations of interleukin-6 and C-reactive protein in patients with benign prostatic disease and prostate cancer is shown in Figure 1P<0.05), interleukin-6 (P<0.01) and C-reactive protein (P<0.01). There was a significant correlation between Gleason score and total PSA , interleukin-6  and C-reactive protein . There was no significant association between total PSA and either interleukin-6 or C-reactive protein.The relationship between Gleason score, PSA, interleukin-6 and C-reactive protein in the patients with prostate cancer is shown in In the present study, there were no significant differences in the circulating concentrations of either interleukin-6 or C-reactive protein between patients with benign prostatic disease and those with prostate cancer. Furthermore, the relationship between interleukin-6 and C-reactive protein was similar in both patient groups.In the cancer patients, there was a significant increase in both interleukin-6 and C-reactive protein concentration with increasing tumour grade. These results are also consistent with previous observations that concentrations of interleukin-6 and its receptor are increased in patients with a Gleason score greater than 7 (An alternative explanation would be that the source of interleukin-6 is the host inflammatory cells. It is therefore of interest that recent studies have shown that a pronounced inflammatory infiltrate in the tumour was associated with poor outcome in patients with prostate cancer (Taken together, these observations confirm that interleukin-6 is predominantly responsible for the elaboration of C-reactive protein. Furthermore, they suggest that interleukin-6 is produced primarily by the host rather than the tumour. Moreover, it may also be that interleukin-6, produced by tumour-infiltrating leucocytes, stimulates tumour cell proliferation and promotes leucocyte recruitment as part of an autocrine growth factor loop .In summary, the results of the present study indicate that the relationship between interleukin-6 and C-reactive protein is similar in patients with benign and malignant prostate disease.


Interleukin-10(IL-10) is a multifunctional cytokine with both immunosuppressive and antiangiogenic functions. Polymorphisms in the IL-10 gene promoter genetically determine interindividual differences in IL-10 production. This study was performed to determined whether polymorphisms in the IL-10 gene promoter were associated with breast cancer in a Chinese Han population.We genotyped 315 patients with breast cancer and 322 healthy control subjects for -1082A/G, -819T/C and -592A/C single nucleotide polymorphisms in the promoter region of the IL-10 gene by polymerase chain reactionerestriction fragment length polymorphism (PCR-RFLP).P = 0.041) and larger tumor size (P = 0.039) at the time of diagnosis. Furthermore, in the haplotype analysis of IL-10 gene, we found that patients carrying ATA haplotype were in higher LN involvement (p = 0.022) and higher tumor stage(p = 0.028) of breast cancer at the time of diagnosis compared with others.There were no significant differences in genotype, allele, or haplotype frequencies in all three loci between patients and healthy controls. Analysis of breast cancer prognostic and predictive factors revealed that the -1082AA genotype was associated with a significantly increased risk of lymph node (LN) involvement (Our findings suggest that IL-10 promoter polymorphisms participate in the progression of breast cancer rather than in its initial development in Chinese Han women. Breast cancer is the most common malignancy threatening the health and life of women and it's incidence has increased in recent years in both developed and developing countries. BiologiIL-10 is an important immunoregulatory cytokine mainly produced by activated T cells, monocytes, B cells and thymocytes. As an immune response modulator, IL-10 can both stimulate and suppress the immune response. NumerouThe gene encoding IL-10 is located on human chromosome 1q31-1q32,9, and iBlood samples were taken from 315 breast cancer cases and 322 non-cancer controls. The case group was recruited between October 2008 and October 2009 from the Shan Dong Cancer Hospital and the PLA 456 Hospital. The patients' pathological and clinical information were obtained from their medical files. All cases were newly diagnosed and previously untreated. The control group consisted of 322 healthy age-matched women who visited the general health check-up division at the two hospitals in the period between October 2008 and October 2009. Selection criteria for controls were no evidence of any personal or family history of cancer or other serious illness. At recruitment, each participant was personally interviewed to obtain detailed information on demographic characteristics and lifetime history of tobacco and alcohol use. All subjects were unrelated ethnic Han Chinese and residents of northern China. The study has been approved by the Institutional Review Boards of Shan Dong Cancer Hospital and the PLA 456 Hospital. Written informed consent was obtained from all participating subjects.2. The PCR profile consisted of an initial melting step of 5 minutes at 94°C, followed by 35 cycles of 30 seconds at 94°C, 45 seconds at 58°C for -1082A/G, 59°C for -819 T/C and 62°C for -592 A/C; 55 s at 72°C; and a final elongation at 72°C for 8 min. The restriction endonucleases MnlI, MaeIII, and RsaI  were used to distinguish the IL-10 gene -1082A/G, -819T/C, -592A/C polymorphisms, respectively  was used for all of the statistical analysis.Genotype and allele frequencies of IL-10 were compared between breast cancer cases and controls by the chi-squared test or Fisher's exact test when necessary. The odds rations (OR) and 95% confidence interval (CI) were calculated to estimate the relative risk conferred by a particular allele and genotype. Demographic and clinical data between groups were compared by chi-squared test and by Student's t-test. Statistical significance was assumed at the P = 0.979 and P = 0.593, respectively), and this suggested that the matching based on these two variables was adequate.The demographics of the cases and controls enrolled in this study are summarized in TableThe genotype and allele frequencies of the IL-10 gene polymorphisms in breast cancer patients and healthy controls are show in TableP = 0.039). In addition, LN involvement was significantly lower in patients harboring at least one G allele at position -1082 A/G (AG and GG genotypes) in comparison to patients with the AA genotype(P = 0.041). . In addition, we also found strong association between tumor size and the ATA haplotypes(P = 0.028). . TableIn this case-control study we evaluated the association between the polymorphisms of the IL-10 promoter and breast cancer in a Han Chinese population. Our data did not show significant differences in allele, genotype and haplotype frequencies between breast cancer patients and healthy controls. In concordance with our study, Howell et al., Smith eIn this study, we found that the frequencies of the -1082 G, -819 C and -592 C alleles among the healthy controls  were similar to those observed in an Asian population,22 but sP = 0.041) and larger tumor size (P = 0.039) at the time of diagnosis. Moreover, in the haplotype analysis of IL-10 gene, we also found that patients carrying ATA haplotype were in higher LN involvement (p = 0.022) and higher tumor stage(p = 0.028) of breast cancer at the time of diagnosis compared with others. The findings suggest that the IL-10 ATA haplotype and -1082AA genotype might be adverse prognostic factors in breast cancer in Chinese Han women.Although we did not find an association of the IL-10 gene polymorphisms with risk of breast cancer, in present study we reported for the first time that the IL-10 promoter haplotypes and -1082 A/G polymorphism were significantly associated with the prognostic and predictive factors of breast cancer in Chinese han women. Our data showed that the -1082AA genotype was associated with a significantly increased risk of lymph node (LN) involvement (IL-10 is a multifunctional cytokine with both immunosuppressive and antiangiogenic functions, which may play varied roles in the pathogenesis and development of breast cancer. Although the genetic control of IL-10 expression is not clearly understood yet, polymorphisms in promoter have been reported to determine interindividual differences in IL-10 production,9. PreviIn conclusion, in this case-control study, we report for the first time that the IL-10 promoter polymorphisms were significantly associated with the prognostic and predictive factors of breast cancer in a Chinese han population. The main finding of our study suggests that IL-10 promoter polymorphisms participate in the progression of breast cancer rather than in its initial development. Considering limited sample size, nonrandom sampling and pitfalls of unknown confounders, further studies with larger sample size from different ethnic origins are required to confirm and extend our observations. In addition, more studies should be carried out to clarify the exact molecular mechanism of IL-10 polymorphisms effects.The authors declare that they have no competing interests.WL, FK and JL designed the study, collected the materials, performed all experiments, YL drafted the manuscript. BS and HW participated in the study and performed the statistical analysis. All authors read and approved the final version manuscript.


In asthma, higher chymase positive mast cell (MC-C) numbers are associated with less airway obstruction. In COPD, the distribution of MC-C and tryptase positive mast cells (MC-T) in central and peripheral airways, and their relation with lung function, is unknown. We compared MC-T and MC-C distributions in COPD and controls without airflow limitation, and determined their relation with lung function.1% predicted 56 [23–75]) and 10 controls were stained for tryptase and chymase. Numbers of MC-T and MC-C were determined in different regions of central and peripheral airways and percentage of degranulation was determined.Lung tissue sections from 19 COPD patients (median [interquartile range] FEV1/VC (Spearman's rho (rs) 0.47, p = 0.05 and rs 0.48, p = 0.05, respectively); MC-C numbers in airway smooth muscle of peripheral airways correlated positively with FEV1% predicted . Both in COPD patients and controls the percentage of degranulated MC-T and MC-C mast cells was higher in peripheral than in central airways , but this was not different between the groups.COPD patients had lower MC-T numbers in the subepithelial area of central airways than controls. In COPD, MC-T numbers in the airway wall and more specifically in the epithelium and subepithelial area of peripheral airways correlated positively with FEVMore MC-T and MC-C in peripheral airways correlate with better lung function in COPD patients. It is yet to determine whether this reflects a protective association of mast cells with COPD pathogenesis, or that other explanations are to be considered. Thirteen out of 19 patients with COPD were treated with oral or inhaled corticosteroids or their combination. Controls without airflow limitation and without corticosteroid use  were former (n = 5) or current (n = 4) smokers; smoking status of 1 control was unknown. Median [range] FEV1% pred of the control group was 103 [90–106], and FEV1/VC ratio 72% [72–75]. Lung function data were missing from one control, a 43-year-old male organ donor, from whom we obtained lung tissue that had not been used for transplantation because of logistic  reasons. An experienced pulmonary pathologist (WT) found no signs of COPD or other significant pathology in lung tissue.Tissue samples were collected from 29 individuals: 19 with COPD and 10 controls without airflow limitation (Table ).Lung tissue samples from GOLD stage IV COPD patients were obtained from tissue remaining after standard pathology protocols in case of lung transplant procedures. The other tissue samples were obtained from subjects undergoing partial resection, (bi)lobectomy, or pneumonectomy for pulmonary carcinoma or metastasis. None of the patients had received chemotherapy. Lung tissue was taken as distant as possible from the tumor, or from non-involved lobes. The study protocol was consistent with national ethical and professional guidelines  alveolar epithelial cells  and chymase  using an immuno-alkaline phosphatase method with goat-anti-mouse immunoglobulin conjugated to alkaline phosphatase as a second step and Fast Red/Napthol AS-Mx as a chromogen providing a bright red reaction product. Double staining was performed to delineate the different compartments. After chymase or tryptase staining, slides were incubated with a mixture of mouse monoclonal antibodies detecting anti-alpha-smooth-muscle actin  and pan-keratin . As a detection step, slides were incubated with rabbit anti-mouse immunoglobulin conjugated to peroxidase. 3'3'di-amino-benzidin together with HWe examined 1 central airway and, depending on the availability, 1 to 5 peripheral airways from each subject. Central airways were defined as cartilaginous airways, maximum from the third generation. Three central airway regions were analyzed separately: 1) epithelium; 2) subepithelial area ; and 3) airway smooth muscle.Peripheral airways were defined as airways with a circumferential size of maximally 6 mm, and a variation of maximal versus minimal diameter ≤ two-fold. Four peripheral airway regions were analyzed separately: 1) epithelium; 2) subepithelial area; 3) airway smooth muscle; and 4) adventitia (area outside smooth muscle to the connection points of the surrounding parenchyma).All areas were measured using the Leica QuantiMed morphometric system  and chymase and tryptase positive, nucleated cells were counted manually within each area and in addition related to all these areas together.The percentage of degranulation was determined for all MC-C and MC-T mast cells, using the method of Carroll et al . In shorA random selection of 10% of all tryptase sections was recounted to determine the intra-class correlation coefficient (ICC), reflecting level of agreement for repeat counts by the same observer. The same was done by two independent observers to determine inter-observer agreement.2 area . In case two or more peripheral airways from one subject were examined, mean cell concentration of all peripheral airways from this subject was used for statistical analysis. Differences between groups were analyzed using Chi-Square or Mann-Whitney U-tests, paired analyses within groups using Wilcoxon-rank-sum-tests, and correlations between cell concentrations and lung function using Spearman's rank correlation coefficient (rs). SPSS 12.0  software was used for statistical analysis. A p-value of 0.05 or less was considered significant.Means and standard deviations or medians with interquartile ranges (IQR) of variables were calculated. Numbers of chymase and tryptase positive mast cells were expressed per mm length of basement membrane (epithelium) or mm1 (% pred) and FEV1/VC values than controls without airflow limitation (both p < 0.001). Gender, smoking status, and pack-years smoking were similar in the two groups (p > 0.05).Patients with COPD had significantly lower FEV2 [0.5–1.2] for the subepithelial area, and 0.6 mm2 [0.4–0.9] for airway smooth muscle.Median [IQR] length of epithelium analyzed in the total study population was 7.5 mm [7.3–7.6], median tissue area 0.7 mm2 [0.02–0.04] for the subepithelial area, 0.02 mm2 [0.01–0.03] for airway smooth muscle, and 0.1 mm2 [0.1–0.2] for the adventitia. Comparable tissue areas were analyzed in the two groups (p > 0.05).The length of epithelium per tissue section analyzed in the total study population was 2.3 mm [2.0–2.8] per peripheral airway. Median tissue area analyzed was 0.02 mmIn central airways, the concentration of MC-T was highest in the subepithelial area in both groups Table . In peri2 [25.7–120.8] versus 108.8/mm2 [83.9–190.2], respectively, p = 0.04). Concentrations of MC-T in any of the other regions of central and peripheral airways were similar in both groups (p > 0.05).The concentration of MC-T was significantly lower in the subepithelial area of central airways from patients with COPD than from controls  without a difference in numbers in smooth muscle.Concentrations of MC-C were highest in the subepithelial area of central airways in both patients with COPD and controls Table . In peri2 versus 0.0/mm2 [0.0–0.3], respectively, p = 0.05). Concentrations of MC-C in any other of the regions of both central and peripheral airways were similar in both groups.The concentration of MC-C was significantly higher in the epithelium of central airways from patients with COPD than from controls, but with very low cell numbers , but not in controls (p = 0.88 and 0.67 respectively).The percentage of degranulated mast cells was higher in peripheral than in central airways for MC-T as well as MC-C (both p < .001), but was not significantly different between COPD and controls.Exclusion of 2 control subjects with missing lung function data did not affect any of the results. A subgroup analysis excluding all COPD patients treated with corticosteroids provided similar results as presented above. The inter-group difference in subepithelial MC-T in central airways was reduced to a trend (p = 0.09), probably due to loss of power.The ICC for the intra-observer variability was 0.96; the inter-observer variability for two independent observers resulted in an ICC of 0.98. This reflects excellent levels of agreement for repeat counts by either the same observer or two independent observers.1/VC . Higher numbers of MC-T in the adventitia of peripheral airways tended to correlate with higher values of FEV1/VC  and FEV1% pred . A higher percentage of MC-T in the total airway wall area of peripheral airways also correlated significantly with higher values of FEV1 and FEV1/VC .Higher numbers of MC-T in the epithelium and in the subepithelial area of peripheral airways correlated significantly with higher values of FEV1% pred or FEV1/VC).No significant correlations existed between the number of MC-T in any of the other airway regions and lung function . Higher numbers of MC-C in the adventitia of peripheral airways tended to correlate positively with FEV1% pred .Higher numbers of MC-C in the smooth airway muscle of peripheral airways correlated significantly with higher values of FEV1% pred or FEV1/VC).We found no significant correlations between the number of MC-C in any of the other airway regions and lung function  in patients with COPD than in controls [1 and FEV1/VC into account, whereas Grashoff's study, conducted before these guidelines were introduced, selected groups solely based on FEV1. Compatible with our findings, no correlation between MC-T in central airways and FEV1 was observed in another study investigating a heterogeneous population of 29 current smokers with and without airflow limitation [Previous studies on the role of mast cells in COPD have not investigated both central and peripheral airways nor the relation of mast cells and their specific granule contents with lung function ,16. We ocontrols . These scontrols ,17 takinmitation , where hmitation . HoweverThere are a few considerations when interpreting our results. Some studies  but not Another point which might have had an impact on our results is that we used lung tissue obtained during surgery for lung cancer. The relation between tumour and mast cells has mainly been studied with respect to mast cells in the tumour tissue and/or effects of mast cells on the tumour . In factHow to put the present results in perspective of the underlying pathophysiology of COPD? The significant relation of MC-T and MC-C with airflow limitation in COPD was present in peripheral and not central airways, compatible with the widely accepted notion that peripheral airways are the major site of airflow limitation in COPD ,22. In lWe realize that there are several arguments against a protective role for mast cells in COPD, since they secrete e.g. proteases, interleukin-8  and tumoIn conclusion, in our view, the strength of the current study is that it is the first to investigate the distribution of tryptase and chymase positive mast cells in central and peripheral airways, and their relation with airflow limitation in patients with COPD. It remains to be clarified whether our findings reflect a beneficial or a detrimental effect of mast cells, or even a combination of both, on lung function in patients with COPD.ASM: alpha-smooth-muscle actin; COPD: Chronic Obstructive Pulmonary Disease; FEV: Forced Expiratory Volume; GOLD: Global Initiative on Lung Disease; IQR: interquartile ranges; MC-C: mast cell-chymase; MC-T: mast cell-tryptase; VC: Vital Capacity.The authors declare that they have no competing interests.MG carried out the data analysis and drafted the manuscript. DP, NHTtH and WT participated in the design of the original study, were responsible for clinical and histological patient data and contributed substantially to the manuscript. BR, ML, MS, and MAL carried out the immunohistology experiments and contributed to the manuscript. JV assisted with data analysis and interpretation, supervised statistics and contributed to the manuscript.


Lactobacillus casei CRL 431 administration improved the resistance to pneumococcal infection in a mouse model.We have previously demonstrated that Lactobacillus casei CRL 431 (L. casei) on the activation of coagulation and fibrinolytic systems as well as their inhibitors during a Streptococcus pneumoniae infection in mice.This study examined the effects of the oral administration of L. casei showed an improvement of alveolo-capillary membrane, lower fibrin(ogen) deposits in lung and decrease in TATc. APTT test and PT, FVII and FX activity were normalized. L. casei group showed lower F levels than control during whole experiment. In the present study no effect of L. casei on the recovery of the inhibitory activity was detected. However, L. casei was effective in reducing PAI-1 levels in BAL and in increasing anti-inflammatory ILs concentration.The alveolo-capillary membrane was damaged and the coagulation system was also activated by the infection. As a consequence, we could see fibrin(ogen) deposits in lung histological slices, increased levels of thrombin-antithrombin complex (TATc) in bronchoalveolar lavage (BAL) and plasma, decrease in prothrombin activity (PT) and prolonged activated partial thromboplastin time test (APTT) values. Factor VII (FVII) and factor X (FX) were decreased in plasma, whereas fibrinogen (F) and factor VIII (FVIII) were increased. The low levels of protein C (PC) in BAL and plasma proved damage on inhibitory activity. The infected animals showed reduced fibrinolytic activity, evidenced by an increase in plasminogen activation inhibitor-1 (PAI-1) in BAL and plasma. The pathogen induced an increase of TNF-α, IL-1β and IL-6 in BAL and serum a few hours after challenge followed by a significant decrease until the end of the assayed period. IL-4 and IL-10 in BAL and serum were also augmented, especially at the end of the experiment. The animals treated with L. casei proved effective to regulate coagulation activation and fibrinolysis inhibition during infection, leading to a decrease in fibrin deposits in lung. This protective effect of L. casei would be mediated by the induction of higher levels of IL-4 and IL-10 which could regulate the anti-inflammatory, procoagulant and antifibrinolytic effects of TNF-α, IL-1β and IL-6. The activation of coagulation and fibrin deposition as a consequence of inflammation is well known, and can be viewed as an essential part of the host defences . The halCurrent evidence from human studies suggests that in lung injury there is augmented tissue factor expression, down regulation of protein C (PC), and higher plasminogen activator inhibitor -1 (PAI-1) levels. Together, these abnormalities shift the intra-alveolar environment from anticoagulant and profibrinolytic to procoagulant and antifibrinolytic .The relationship between inflammation and the coagulation system is a process in which inflammation leads not only to the activation of coagulation, but coagulation also considerably affects inflammatory activity. Besides, an insufficiently controlled response can lead to a situation in which coagulation and thrombosis contribute to disease .Hence, modulation of fibrin deposition through coagulation and fibrinolysis regulation may be an important therapeutic target.Lactobacillus casei CRL 431 to mice infected intranasally with Streptococcus pneumoniae (S. pneumoniae) facilitated clearance of the pathogen and modulated the inflammatory immune response with less damage to lung tissue [Probiotic lactic acid bacteria have several inmunomodulatory effects ,6 and ang tissue .Lactobacillus casei CRL 431 on the activation of coagulation during a S. pneumoniae infection in a mouse experimental model.Considering the relevant participation of the relationship inflammation-coagulation in the severity of pneumococcal pneumonia , the preLactobacillus casei CRL 431 (L. casei) was obtained from the CERELA culture collection. It was cultured for 8 h at 37°C  in Man-Rogosa-Sharpe broth , and the bacteria were harvested through centrifugation at 5,000 rpm for 10 min and then washed three times with sterile 0.01 M phosphate buffer saline (PBS), pH 7.2 [, pH 7.2 .Capsulated pneumococcus (serotype 14) was isolated from the respiratory tract of a patient from the Department of Clinical Bacteriology of the Niño Jesús Children's Hospital in San Miguel de Tucumán, Argentina. Pneumococci serotypification was performed in Administración Nacional de Laboratorios e Institutos de Salud-ANLIS "Dr. Malbran", Buenos Aires, Argentina.Six-week-old Swiss albino mice were obtained from the closed colony kept at CERELA. They were housed in plastic cages at room temperature. Each assay was performed in groups consisting of 25-30 mice (5-6 for each day before and after infection). The Ethical Committee for Animal Care at CERELA and Universidad Nacional de Tucumán approved the experiments.L. casei was administrated for 2 consecutive days at a dose of 109 cell/mouse/day [L. casei was suspended in 5 ml of sterile 10% non-fat milk (NFM) and added to the drinking water (20% v/v). The control group received sterile NFM in the same conditions as the test group. All mice were fed a conventional balanced diet ad libitum.S. pneumoniae was grown according to Racedo et al. [6 CFU (log-phase) of S. pneumoniae in PBS into each nostril and allowing it to be inhaled. To facilitate migration of the inoculum to the alveoli, mice were held in a head-up vertical position for 2 min. Animals were sacrificed on day 0 (before infection) and at different times post-infection.o et al. . At the L. casei for 2d showed a faster clearence of the S. pneumoniae [The pathogen was detected in lung and blood samples of control mice throughout the period assayed, while the group fed with eumoniae . After t2O2 to inhibit the activity of endogenous peroxidase in the lungs . The sections were incubated in 10% normal sheep serum and then exposed to sheep antimouse fibrinogen . After washes, slides were incubated with donkey antisheep IgG peroxidase conjugate (Sigma-Aldrich Co). Peroxidase activity was detected with a 3,3'-diaminobenzidine peroxidase substrate solution (Sigma-Aldrich Co), after which a light counterstain with hematoxylin was performed [Fibrin(ogen) deposition in pulmonary tissue was determined by immunohistochemical techniques. Lung samples from both groups were fixed in 4% (v/v) formalin saline solution, dehydrated, embedded in Histowax  and cut into 4 μm serial sections. For fibrin(ogen) immunostaining, lung sections were deparaffinized and endogenous peroxidase activity was quenched with a solution of methanol/0.03% Herformed .BAL samples were obtained according to the technique described previously . BrieflyA measure to quantitate increased permeability of the bronchoalveolar-capillarity barrier was determined colorimetrically based on albumin binding to bromocresol green using an albumin BCG diagnostic kit . The results were expressed as mg/mL.An indicator of general cytotoxicity was determined by measuring the formation of a reduced form of nicotinamide adenine dinucleotide using Roche Diagnostic reagents and procedures . The results were expressed as U/L of BAL fluid.et al [Blood samples were obtained through cardiac puncture and were collected in a 3.2% solution of trisodium citrate at a ratio of 9:1. Plasma was obtained according to Agüero et al . Prothroet al .® protein C, Chromogenix, Mölndal, Sweden; STACHROM® PAI, Diagnostica Stago, Asnières, France). TATc, PC and PAI-1 levels were measured in BAL and plasma samples.Thrombin-antithrombin complexes (TATc), markers of coagulation system activation, were determined by enzyme-linked immunosorbent assay (ELISA) technique according to manufacturer's instructions . PC and PAI-1 activities were measured by chromogenic substrate assays , interleukin-1β (IL-1β), IL-4, IL-6 and IL-10 concentrations were measured with commercially available ELISA kits according to the manufacturer's recommendations .P < 0.05.Experiments were performed in triplicate  and results were expressed as means ± SD. After verification of a normal distribution of data, 2-way ANOVA was used. Tukey's test (for pairwise comparisons of the means) was used to test for differences between the groups. Differences were considered significant at S. pneumoniae caused increases in BAL albumin concentration and LDH activity in both groups, but these parameters were significantly lower in L. casei treated mice  deposits in the pleura. These deposits reached their highest intensity at 10d post-infection Figure . In the L casei showed fibrin deposits of only in the pleura, with a focal pattern and lower intensity than in the C group.The animals treated with L. casei, which remained within the normal range since d 2 post-infection.TATc levels were increased in BAL from both experimental groups, showing highest values on d 1 post-infection Figure . Then, TL. casei, returned to normal values on d 5 after challenge, whereas the control group continued with higher values  than control mice until the end of the experiment  than in the control (d 10).Levels of PAI-1 in BAL increased after infection in both experimental groups, reaching a maximum on d 1 post challenge Figure . HoweverSystemic PAI-1 levels showed a similar decrease in both groups, reaching normal values on d 10 post-infection Figure .L. casei group. Afterwards, the values of TNF-α and IL-1β decreased gradually until they returned to base levels on d 5, whereas IL-6 concentration remained elevated with significantly higher values in the control group  to declare in relation to this manuscript.CH did the experimental work, the data analysis and prepared the manuscript; JV contributed to the drafting of the paper; HZ contributed with the experimental work; SA contributed with the designs of study; GA revised the manuscript for the intellectual content and gave final approval. All authours have read and approved the final version of the manuscript.


P<0.001).The value of combining Dukes' stage and C-reactive protein to form a cumulative prognostic score was assessed in 147 patients undergoing potentially curative resection for colorectal cancer. The cancer-specific survival rates at 3 years for patients with a cumulative prognostic score of 0, 1 and 2 were 100, 77 and 40%, respectively (HR 4.76, 2.78–8.15,  Colorectal cancer is the second commonest cause of cancer death in North America and Western Europe. Each year in the UK, there are approximately 27 000 new cases and approximately 18 000 deaths attributable to the disease. Overall survival is poor; even in those who undergo potentially curative resection, only half survive 5 years . Grouping of the variables age, albumin and C-reactive protein was carried out using standard thresholds .Survival (cancer-specific) analysis was performed using the Cox proportional hazard model. Deaths up to September 2003 have been included in the analysis. Multivariate survival analysis was performed using a stepwise backward procedure to derive a final model of the variables that had a significant independent relationship with survival. To remove a variable from the model, the corresponding n=147) who underwent potentially curative surgery for colorectal cancer are shown in Table 1The baseline clinicopathological characteristics of the patients , sex (P<0.05), Dukes' stage (P<0.001), elevated circulating C-reactive protein concentrations (P<0.001) and venous invasion (P<0.05) were associated with poor cancer-specific survival. On multivariate analysis, including the above variables, age , Dukes stage  and C-reactive protein  were significantly associated with cancer-specific survival.On univariate analysis, increased age (−1.Since the magnitude of the covariates of Dukes' stage (1.70) and C-reactive protein (1.45) were similar, this indicates that a one unit increase in C-reactive protein had approximately the same relative risk as a one unit increase in pathological stage and that they could be simply added to form a prognostic score. Such a cumulative prognostic score was therefore constructed by assigning one point for each of the following criteria: Dukes' stage C and C-reactive protein >10 mg lP<0.001).The relationship between stage, C-reactive protein concentration, the cumulative prognostic score and cancer-specific mortality is shown in Several studies have shown that elevated circulating C-reactive protein concentrations are associated with poor survival in patients with colorectal cancer . HoweverIn the present study, both Dukes' stage and C-reactive protein concentrations were independently associated with cancer-specific survival. These results are consistent with those of Nielsen and co-workers (2000) who, in a cohort of almost 400 patients undergoing resection for Dukes B and C tumours, also demonstrated that C-reactive protein was a Dukes' stage independent prognostic factor. The mechanism by which a systemic inflammatory response might influence cancer survival is not clear. However, it is known that as part of the systemic inflammatory response, there is a release of pro-inflammatory cytokines and growth factors which may promote tumour growth .In the present study, when C-reactive protein concentrations were combined with Dukes' stage to form a new prognostic score, the combined score improved the prediction of cancer-specific survival. The addition of C-reactive protein differentiated between low- and high-risk Dukes' B and low- and high-risk Dukes' C patients. Cancer-specific survival at 3 years ranged from 100% in patients with Dukes' B tumours and a normal C-reactive protein concentration to 40% in patients with Dukes' C tumours and an elevated C-reactive protein concentration. Cancer-specific survival in patients with Dukes' B tumours and an elevated C-reactive protein concentration was similar to that of patients with Dukes' C tumours and a normal C-reactive protein concentration. Therefore, this cumulative prognostic score may be useful in identifying high-risk Dukes' B patients suitable for adjuvant therapy.The results of the present study indicate that this simple prognostic score, which reflects both the contribution of the tumour and the host response, differentiates between low- and high-risk Dukes' B and C tumours in patients undergoing potentially curative resection for colorectal cancer.


Some clinical studies indicate that ECP gathers in the bronchial tract of asthma sufferers, damages bronchial and airway epithelial cells, and leads to in breathing tract inflammation; therefore, prevention of the cytotoxicity caused by ECP may serve as an approach to treat airway inflammation. To achieve the purpose, reduction of the ECP-cell interactions is rational. In this work, the Chinese herbal combinative network was generated to predict and identify the functional herbs from the pools of prescriptions. It was useful to select the node herbs and to demonstrate the relative binding ability between ECP and Beas-2B cells with or withour herb treatments.The eosinophil cationic protein (ECP) is cytotoxic to bacteria, viruses, parasites and mammalian cells. Cells are damaged Eighty three Chinese herbs and prescriptions were tested and five effective herbs and six prescription candidates were selected. On the basis of effective single-herbal drugs and prescriptions, a combinative network was generated. We found that a single herb, Gan-cao, served as a node connecting five prescriptions. In addition, Sheng-di-huang, Dang-guei and Mu-tong also appeared in five, four and three kinds of prescriptions, respectively. The extracts of these three herbs indeed effectively inhibited the interactions between ECP and Beas-2B cells. According to the Chinese herbal combinative network, eight of the effective herbal extracts showed inhibitory effects for ECP internalizing into Beas-2B cells. The major components of Gang-cao and Sheng-di-huang, glycyrrhizic acid and verbascose, respectively, reduced the binding affinity between ECP and cells effectively.Since these Chinese herbs reduced the binding affinity between ECP and cells and inhibited subsequent ECP entrance into cells, they were potential for mitigating the airway inflammation symptoms. Additionally, we mentioned a new concept to study the Chinese herbs using combinative network in the field of systems biology. The functional single herbs could be identified from the set of prescriptions. N-terminus of ECP is bactericidal and disrupts lipid membrane of the target cells [Eosinophil cationic protein (ECP) is a human protein secreted from activated eosinophils and released into biological fluid under inflammatory conditions . It beloet cells .via processes of pore formation, permeability alterations and membrane leaking [Some hypothesized mechanisms of ECP-triggered cell damage suggested that ECP destabilized cell membrane  leaking . ECP con leaking  or the i leaking . In clin leaking . The ECP leaking  and cyst leaking . During  leaking . It was  leaking . We have leaking  and bron leaking , respect leaking . in vitro [Since cell damage caused by ECP is one of the significant features of airway inflammation, it is of great interest in discovering practical substance to neutralize the cytotoxicity of ECP. As is known, the cytotoxicity of ECP can be blocked by specific antibodies and heparin in vitro ,24, the in vitro -27. For a long time, Chinese people have used many herbs and prescriptions to treat asthma and airway inflammation. In Chinese medical science, several prescriptions have been available for supportive care against the diseases caused by the patients’ own immune system. Asthma sufferers can be treated by supporting approaches to regulate the whole body’s immunity, but these strategies are too slow to respond effectively in the emergent events. In this report, we characterized some of the Chinese herbs which inhibited interactions between ECP and the airway epithelial cell line so that the cytotoxicity of ECP might be reduced or eliminated. According to the wet lab experimental data, we generated Chinese herbal combinative network to identify the node herbs which were the components of effective prescriptions. In total 83 kinds of Chinese prescriptions and herbs recommended by Chinese medical doctors were screened using ELISA method. The decrease in binding affinity between ECP and bronchial epithelial cells could be quantified. Moreover, the effective Chinese herbs also inhibited the ECP translocating into Beas-2B cells at very low concentration. Most of the major compounds of effective Chinese herbs also showed inhibitory effects on reducing the binding affinity between ECP and cells. They may serve as the potent drugs for the development of anti-asthma or anti-inflammation treatments.    In the first step, thousands of Chinese herbs and prescriptions were prescreened by Chinese medical doctors. After prescreening, two herbal sets containing 90 and 83 kinds of herbs or prescriptions were selected and purchased from Sun Ten Pharmaceutical Co. Ltd. and Chuang Song-Zong Pharmaceutical Co. Ltd., separately. Because of the manufacture process, the single-herbal drugs and prescriptions from Chuang Song-Zong Company were boiled, cooled and concentrated to viscous formulation; however, the drugs from Sun Ten Company appeared clear. After ELISA analysis, the drugs from Sun Ten Company showed less effect on inhibiting the ECP-cell interactions, possibly due to the difference between the two manufacture processes. As a result, the herbs produced by Chuang Song-Zong Company were screened to select the effective candidates for reducing the binding affinity of ECP and cells. The English and Chinese names of herbs used in this study was listed in Additional file via blockage of the ECP-cell interactions. ECP was tagged with MBP for antibody probing in the ELISA process. As shown in Fig. Fig. Fig. Fig. Fig. Fig. Fig. The cell-based ELISA platform is employed to screen for drugs that can reduce or inhibit the interactions between ECP and Beas-2B cells. It is expected that the cytotoxicity of ECP can be inhibited Fig. Fig. ; Goat anti-mouse HRP-from Jackson (USA); Goat anti-mouse FITC-conjugated from Abcam (USA); Hoechst 33342 from Molecular Probes (USA); RPMI 1640 medium from Sigma-Aldrich (USA); fetal bovine serum (FBS) from Gibco/Invitrogen (USA) and tetramethylbenzidine (TMB) from KPL (USA).2 and 95% air.Beas-2B, a human bronchial epithelial cell line (ATCC CRL-9609), was cultured in RPMI-1640 medium supplemented with 10% heat-inactivated fetal bovine serum at 37°C in an incubator containing 5% COE. coli BL 21(DE3) pLysS  for protein expression. Twenty milliliters of the culture grown overnight were inoculated into 500 ml of LB media containing 100 µg/ml carbenicillin  and 50 µg/ml chloramlphenicol , followed by incubation at 37°C for 6 hr till the OD600 reacβ-D-1-thiogalactopyranoside (IPTG)  was added to a final concentration of 1 mM, and the bacteria were harvested after a 4-hr induction at 30°C. Recombinant MBP-ECP was fractioned in the soluble portion of bacterial lysate. MBP-ECP was purified using amylose affinity chromatography  and switched to PBS buffer. To remove the impurity, heparin affinity chromatography was utilized. Then, MBP-ECP was eluted using 100 mM phosphate buffer (pH 7.4) containing 1 M NaCl. The purified MBP-ECP was concentrated and the buffer was changed to PBS by Amicon Ultra-15. The protein concentration of MBP-ECP was determined by BCA assay kit  and the MBP was also purified and served as the negative control for cell-based ELISA.The pMAL-c2X-ECP was transformed into g at 4°C for 20 min, and the supernatant was transferred to a new tube and stored at -20°C. In the cytotoxcity assay, the herbs were filtered by 0.22 μm.The extracts of Chinese herbs were viscous, so they were diluted into distilled water to a final concentration of 10% (w/v). Afterwards, the mixture was centrifuged at 12,000 4Cl and reblocked with 2% BSA/PBS, the MBP-ECP was probed by mouse anti-MBP mAb at 25°C for 1 hr and anti-mouse HRP-conjugated secondary antibody was used to probe the anti-MBP mAb. One hundred microliters of TMB were added into each well prior to incubation for 10 min in the dark. After terminating the reaction by adding 100 μl 2N HCl, the absorbance for 450 nm was measured using ELSA reader . The 450A value indicating the interaction between MBP-ECP and cells was set as 100% and the relative fold of ECP-cell binding reduction was calculated using Microsoft Office Excel 2007.The binding affinity between MBP-ECP and Beas-2B cells in the presence of various Chinese herbs was measured using cell-based ELISA. One microliter of the processed herbs was diluted into 49 μl serum-free RPMI-1640 medium containing 100 nM MBP-ECP. The Beas-2B cells with 70% confluent monolayer in each well of a 96-well plate  were washed with PBS and blocked with 2% BSA/PBS at 4°C for 1 hr. Next, the cells were incubated in 50 μl serum-free media and mixed with 50 μl herb/MBP-ECP mixture at 4°C for 1 hr. After being washed with PBS, fixed with 2% PFA/PBS, quenched with 50 mM NHTable Beas-2B cells were seeded on coverslips into 24-well culture plates  in RPMI 1640 medium containing 10% FBS and incubated for 2 days to form a monolayer at the bottom of plates. In addition, cells were incubated at 37°C for 1 day in serum-free RPMI 1640 medium for serum starvation. Eight herbal extracts, Tao-ren, Sie-cao, Wei-ling-sian, Gan-cao, Jhih-mu, Dang-guei, Sheng-di-huang, Mu-tong, at concentrations of 0.1% or 0.01% (w/v) were mixed with 200 nM MBP-ECP for 30 min at room temperature and treated with Beas-2B cells at 37°C for 1 hr. The major component of Chinese herbs listed in A combinative network was generated according to the ELISA screening results. Each herb of the prescriptions was listed and boxed. The network was generated using CellDesigner 4.0.1  and modi4Cl in PBS for 10 min, and then permeabilized with 0.5% (v/v) Triton-X-100 for 5 min at room temperature in turn. The cells were blocked with 2% BSA in PBS for 30 min at room temperature, and labeled with primary antibodies diluted in 0.5% BSA in PBS for 1 hr at 25°C, followed by washing two times with 0.05% (v/v) Triton-X-100 and one time with PBS, and then incubated with secondary antibodies diluted in 0.5% BSA in PBS for 1 h at 25°C. Nuclei were stained with Hoechst 33342 during the last 5 min of incubation. Cells were then washed two times with 0.05% Triton-X-100 and one time with PBS, and coverslips were mounted by ProLong® Gold antifade reagent (Invetrogen). Confocal microscopy was performed employing an Axiovert (Zeiss) confocal microscope using the 63 × Plan-NEOFLUAR oil immersion objective. After drug treatment, cells were extensively washed with PBS, fixed with 2% paraformaldehyde (PFA) in PBS for 15 min, quenched with 50 mM NHECP: eosinophil cationic protein; MBP: maltose-binding protein; RNase: ribonuclease; ELISA: enzyme-linked immunosorbent assayThe authors declare that they have no competing interests.HTC conceived this study and drafted the manuscript. LJT conducted the cell-based ELISA and drug screening. TJH finished the confocal microscope studies. BTK participated in the drug screening. WYL helped to pre-screen the Chinese herbs in the initial design. TCF and MDTC helped the establishment of the cell-based ELISA. TWP participated in the network generation.The English and Chinese names of herbs used in this study.Click here for fileRelative binding affinity between ECP and cells is decreased by adding different concentrations of the major components of the Chinese herbs. Each experiment is duplicated.Click here for file


With technological advances in high-throughput genotyping, it is not unusual to perform hundreds of thousands of tests for each phenotype. Thus, correction to control type I error is essential. The false-discovery rate (FDR) has been successfully used in genome-wide expression data. However, its performance has not been evaluated for association analysis. Our objective was to analyze the Genetic Analysis Workshop 15 simulated data set, with answers, to evaluate FDR for genome-wide association and fine mapping. In genome-wide analysis, FDR performed well, with good localization of positive results. However, in fine mapping, all tested methods performed poorly, producing a high proportion of significant results. Thus, caution should be used when employing FDR for fine mapping. Multiple testing within research studies takes many forms, including multiple comparisons regarding a single outcome, multiple outcomes, or multiple secondary analyses. Each scenario increases the overall probability of false-positive outcomes. With technological advances in high-throughput genotyping, it is not unusual to perform hundreds of thousands of tests for each phenotype in genome-wide association studies. Thus, strategies are required to control type I error rate.q-value, the FDR analogue of the p-value [q-value is the proportion of significant single-nucleotide polymorphisms (SNPs) that will be false positives for a given threshold. For example, if we want to review SNPs with q-values ≤ 0.05 and this yields 200 SNPs, then approximately 10 of these 'significant' SNPs are expected to be null. Storey assumes that null p-values are uniformly distributed, and can thus be used to determine the q-value. These assumptions are valid for genome-wide expression data [The false-discovery rate (FDR) was originally proposed by Benjamini and Hochberg as a method to adjust for multiple comparisons . Less co p-value . The q-vp-values and FDR performance has not been explored. Therefore, our objective was to analyze the Genetic Analysis Workshop 15 (GAW15) simulated data set, with answers, to evaluate FDR for genome-wide association and fine mapping.In genome-wide linkage and association studies, it is common practice to follow up previously identified linkage and association signals with dense genotyping in that area . Genotyping a large number of SNPs in a region may introduce non-independence due to linkage disequilibrium (LD). Additionally, we would expect an increase in the number of true positives, as these regions were selected because of previous evidence of linkage or association. Unfortunately, the impact of non-independence and increased true positives on the distribution of null GAW15 simulated rheumatoid arthritis (RA) data consists of 1500 families with an affected sibling pair and 2000 unaffected controls. The authors had knowledge of the simulated answers at the time of analyses.All 9187 SNPs on 22 autosomes were analyzed in 11 replicates (90–100). The dense map of 17,820 SNPs on chromosome 6 was used to further model an analysis of follow-up of a known linkage peak region. Of these SNPs, 2094 within 10 cM flanking the causative allele at locus DR, as provided in the answers, were selected for analysis.p-values for each replicate were imported into QVALUE and analyzed using default settings and a FDR of 0.05 [0 (the probability that a given hypothesis is truly null). To compare methods, FDR was also estimated using the method of Benjamini and Hochberg [p-values [p-value threshold of 0.05 and Bonferroni corrected thresholds of 0.05/9187 = 5*10-6 for genome-wide and 0.05/2094 = 2*10-5 for fine mapping.CCREL in the computer program R  was used of 0.05 . To compHochberg  and a rop-values . Resultsp-values, we inspected the p-value distribution from all replicates combined. To explore the empirical false-positive (FP) rate, SNPs identified as significant in the genome-wide association analysis were examined for proximity to any of the modeled causative loci. Significant SNPs within either 10 cM or 5 cM of causative loci were labeled 'true positives' (TP), and the proportion of all significant SNPs labeled 'true positive' and 'false positive' (where FP = 1 - TP) was calculated.To examine the underlying assumption of a uniform distribution of r2. Spearman's rank correlation between LD and p-values for the association was used to determine if the p-values were related to LD.Because non-independence could also influence the results, we examined LD between the SNP at the DR locus and 186 SNPs flanking DR. LD was calculated using p ≤ 0.05 compared to 0.27% and 0.21% of SNPs using FDR and Bonferroni adjustment, respectively. Estimated π0 averaged 0.99  were >10 cM and 5003 (90%) were >5 cM from a causative locus. Even on chromosomes with a causative locus, the 5-cM FP rate was marginally lower on only three of the six chromosomes . Using FDR thresholds, only 9 of the 272 SNPs were >5 cM from a causative locus, for a 5-cM FP rate of 3.3%, with the vast majority (96.3%) within 3 cM of a causative locus. Bonferroni thresholds identified 217 significant SNPs, all but one of which (0.46%) was within 3 cM of causative loci. Loci that mediated RA hazard or increased RA severity were not identified using any method.FP rates for significant SNPs differed greatly by method. For the nominal p ≤ 0.05 compared with 46.4% and 21.4% of SNPs using FDR and Bonferroni, respectively. A cluster of 186 significant SNPs spanned 2 Mb around DR, but these accounted for less than 20% of significant associations.In contrast, using the dense chromosome 6 SNPs, an average of 45.2% of SNPs was significant at 0 was much lower, averaging 0.41  was 0.20, with very little correlation between LD and the p-value of the association .In the SNPs flanking DR, average LD , and similar to Bonferroni adjusted p-value thresholds. Defining false positives as greater than 5 cM from a causal locus, SNPs identified by FDR exhibited only a 3.3% FP rate; thus, FDR was empirically more conservative than expected in this analysis.Genome-wide FDR reduced the number of results identified as significant compared with unadjusted p-value methods all produced a high proportion of significant results. This breakdown in FDR in fine mapping may be due to the extremely skewed distribution of p-values seen in the region harboring disease-causing alleles. Pounds and Cheng [p-values in FDR estimation. When FDR was estimated using a robust methodology, there was a marked reduction in the number of significant tests; however, a large number of significant results remained. Likewise, the low correlation between pair-wise LD and p-values suggests that the large number of significant findings was not due to LD. However, because the causative locus was modeled as a tri-allelic marker, the pairwise LD between bi-allelic SNPs may not capture the true extent of LD with the DR locus.For fine-mapped regions, FDR and nd Cheng  have notp < 10-300) between the DR locus  and RA. Because GAW15 Problem 3 was modeled on results from RA analyses, it is possible that other diseases may exhibit similarly strong associations that persist over a large number of SNPs. This may be especially true when causal variants are multi-allelic, such as the relationship between Kringle repeat number in the apolipoprotein A gene and serum Lp(a) [We suspect that these findings were due to the very strong association  declare that they have no competing interests.